Efficacy of Centanafadine for Executive Functioning and Learning Problems in Children and Adolescents with Attention-Deficit/Hyperactivity Disorder

Abstract

Objective:

The objective of these analyses was to evaluate the efficacy of centanafadine for treatment of attention-deficit/hyperactivity disorder (ADHD) associated features of executive functioning and learning problems measured over 6 weeks in children and adolescents with ADHD.

Methods:

Two phase 3, randomized, double-blind, placebo-controlled trials were conducted in children and adolescents with a primary diagnosis of ADHD at sites in the United States and Canada. Data presented here are from the total population of children aged 6–12 years (NCT05428033) or adolescents aged 13–17 years (NCT05257265) who received high-dose centanafadine or placebo for 6 weeks. Secondary and other efficacy endpoints assessed change from baseline in the Conners 3–Parent Short Executive Functioning and Learning Problems Content Scale T-scores and the Conners 3–Self-report Short Learning Problems Content Scale T-scores, all analyzed using a mixed-effect model for repeated measures. Clinically meaningful within-patient change and findings from a caregiver and/or adolescent self-report exit survey are also presented.

Results:

Overall, 76.5% (367/480) of children (mean age 9.2 years, 41.7% female) and 80.8% (371/459) of adolescents (mean age 14.7 years, 40.7% female) completed their respective studies. There were clinically significant improvements with centanafadine treatment observed as early as Week 1 in the Conners 3–Parent Short Executive Functioning (T-score least squares mean change from baseline [standard error] to Week 6: children, −6.8 [1.1] vs. −11.3 [1.1], p = 0.0026 and adolescents, −8.1 [1.0] vs. −13.0 [1.0], p = 0.0003) and Learning Problems content scale T-scores when compared to placebo (children, −8.2 [1.0] vs. −2.8 [0.9], p < 0.0001 and adolescents, −8.0 [0.9] vs. −3.1 [0.9], p < 0.0001). Similar changes from baseline were observed by adolescent self-report (−6.1 [0.9] vs. −2.5 [0.9], p = 0.0023). Compared to placebo, there was a 50% and 88% greater chance of experiencing clinically meaningful within-patient change in executive functioning for children and adolescents, respectively, and a 79% and 81% chance for experiencing clinically meaningful within-patient change in learning problems, respectively. Exit survey data support findings from clinical outcome measures.

Conclusions:

In addition to its impact on the core symptoms of ADHD, centanafadine improved executive functioning, learning problems, and impact on daily tasks in children and adolescents with ADHD.

Keywords

attention-deficit/hyperactivity disorder centanafadine children adolescents executive functioning learning problems

Introduction

Attention-deficit/hyperactivity disorder (ADHD) is a chronic neurodevelopmental disorder that is common among both pediatric and adult populations (Faraone et al., 2024) and impacts daily functioning and quality of life (Faraone et al., 2021a; Varrasi et al., 2022). In addition to the core ADHD behavioral symptoms (inattention, impulsivity, and hyperactivity) (Faraone et al., 2024), individuals living with ADHD can experience ADHD associated features, including executive functioning deficits and emotional dysregulation.

Executive functioning refers to a set of cognitive processes that include working memory (temporarily storing and using information to complete a task), inhibitory control (resisting impulses and distractions), cognitive flexibility (adapting to changing situations or perspectives), planning, shifting, and organization (Rabinovici et al., 2015; Wilens et al., 2024; Willcutt et al., 2005). Executive dysfunction is also associated with emotional dysregulation, as executive control processes such as inhibitory control and working memory are central to the regulation of emotional responses (Groves et al., 2022). Executive dysfunction is common in ADHD; for example, in one study, more than 89% of children with ADHD were found to manifest some impairment in executive functioning (Kofler et al., 2019). Data also suggest that approximately one-third of children and adolescents with ADHD manifest clinically significant executive dysfunction (Biederman et al., 2004; Wilens et al., 2024).

Executive functioning is fundamental to learning as it provides the cognitive foundation necessary for success in a traditional academic setting. Executive dysfunction can lead to loss of focus, distractions, and an inability to shift between tasks (Rabinovici et al., 2015; Wilens et al., 2024; Willcutt et al., 2005). Planning and organization skills help manage time, materials, sequencing, and assignments, while inhibitory control allows students to resist impulses, including physical, verbal, and emotional impulses, as well as staying on task (Rabinovici et al., 2015). Specifically, executive functioning has been shown to be important for both reading comprehension (Avramovich and Yeari, 2024; Miranda-Casas et al., 2010) and mathematical problem-solving (Kaskens et al., 2022; Viterbori et al., 2017). Deficits in cognitive-executive-functioning processes may underlie learning difficulties, with variable effects across individuals (Avramovich and Yeari, 2024; Soto et al., 2021; Tamm et al., 2021).

Executive functioning can be measured using clinical and neuropsychological scales, testing, or a combination of both. Clinical scales (e.g., Conners 3 [Conners, 2008], Behavior Rating Inventory of Executive Function [Gioia et al., 2000]) usually involve questionnaires or interviews that rate behaviors related to executive functioning in daily life. These can be completed by investigators/clinicians, caregivers, or participants. In contrast, neuropsychological scales (e.g., Cambridge Neuropsychological Test Automated Battery; Cambridge Cognition, 2019) consist of structured cognitive tasks designed to objectively assess specific executive processes. Standardized rating scales and neuropsychological testing may identify different individuals as having executive dysfunction and may also capture variability in the specific components of executive functioning affected within the same individual, highlighting the heterogeneity of this ADHD associated feature (Snyder et al., 2015; Soto et al., 2020).

Stimulants are among the first-line therapies used to treat ADHD (Childress, 2022; Cortese, 2023). Although these treatments are effective for treating core ADHD symptoms, prior work indicates that ∼25% of individuals who responded to methylphenidate continued to exhibit executive function impairments, suggesting incomplete improvement and the potential for ongoing functional impairment despite treatment (Biederman et al., 2011). Compared with the large effect sizes observed for core ADHD symptoms, analyses suggest that small-to-moderate effect sizes are observed for executive functioning (Isfandnia et al., 2024). Similarly, nonstimulants were also shown to have variable effects on executive functioning. For example, atomoxetine has been shown to improve nonverbal executive functioning (Brown et al., 2011; Gau and Shang, 2010) in children and adults with ADHD, although it appears to have a lower impact on response selection/inhibition (Wu et al., 2021) compared with methylphenidate, and no impact on lexical decision (deciding if a string of letters is a word or not) (de Jong et al., 2009). Moreover, in a post hoc analysis of four randomized clinical trials with viloxazine extended-release, 39% of individuals were considered executive function treatment responders with a small effect size of 0.24 (Faraone et al., 2021b).

In terms of effects on learning abilities, studies assessing the impact of available ADHD treatments have failed to show clinically relevant improvements, with children with ADHD consistently showing academic deficits compared with their peers (Advokat and Scheithauer, 2013; Varnet Perez et al., 2025). Methylphenidate has been shown to improve classroom productivity, but these benefits failed to translate into academic learning (Pelham et al., 2022; van der Schans et al., 2017). Since the nature and severity of executive dysfunction vary widely among individuals living with ADHD, currently available treatments may not be equally effective, particularly when specific components of executive functioning are differentially impaired, highlighting a significant unmet need. Developing new therapies that can more comprehensively address both the core symptoms of ADHD and associated features such as executive dysfunction is essential.

Monoamine neurotransmitters (dopamine, norepinephrine, and serotonin) have been shown to modulate activity patterns in the key brain regions involved in ADHD (Azizi, 2022; Faraone et al., 2024), and alterations in their signaling may mediate the varied symptoms and multiple comorbidities associated with ADHD (Jackson et al., 2025; Kanarik et al., 2022; Zhang et al., 2023). All three monoamines have been implicated in some aspects of executive functioning (Aznar and Hervig, 2016; Kanarik et al., 2022; Klaus and Pennington, 2019); however, many of the current treatments for ADHD mostly impact norepinephrine and/or dopamine signaling (Pliszka, 2005).

Centanafadine is a norepinephrine, dopamine, and serotonin reuptake inhibitor (NDSRI) in development for the treatment of ADHD. It has been shown to be efficacious, with a favorable safety profile and well-tolerated in phase 2 and phase 3 clinical trials for ADHD in children, adolescents, and adults (Adler et al., 2022; Ward et al., 2025a, 2025b; Wigal et al., 2020). Given the importance of addressing executive functioning in ADHD, the objective of these analyses was to evaluate the efficacy of centanafadine for the treatment of ADHD associated features of executive functioning and learning problems from the total population of children and adolescents from two phase 3 clinical trials of centanafadine.

Materials and Methods

Trial design

This publication of prespecified analyses, using secondary and other efficacy outcome data, was derived from two phase 3, randomized, double-blind, placebo-controlled trials conducted in children and adolescents with ADHD. The details of both trials have been previously published (Ward et al., 2025a, 2025b). In brief, a trial in children aged 4–12 years (registered at ClinicalTrials.gov; NCT05428033) was conducted at 58 sites in the United States and Canada. At the time of writing, data analysis for the younger cohort (4–5 years of age) is ongoing, so only data from the older cohort (6–12 years of age) are reported here. A similar trial was conducted in an adolescent population, with participants aged 13–17 years (registered at ClinicalTrials.gov; NCT05257265) at 48 sites in the United States and Canada.

Both trials had a duration of ∼11 weeks, which included a screening period (≤4 weeks, including a washout period for participants taking stimulant ADHD medications at screening), a double-blind treatment period (6 weeks), and a follow-up period (7 days [+2]). The outcomes reported herein are based on the 6-week double-blind trial. Both trials were completed in compliance with the study protocol, Good Clinical Practice Guidelines, the principles derived from the Declaration of Helsinki, and all other applicable local laws and regulations. All trial documentation was reviewed and approved by an Institutional Review Board (Advarra Central IRB) or Independent Ethics Committee for each investigational site/country.

Participants

A complete list of inclusion and exclusion criteria has been previously published (Ward et al., 2025a, 2025b). In brief, key eligibility criteria for both trials were a primary diagnosis of ADHD based on Diagnostic and Statistical Manual of Mental Disorders, Fifth Edition (DSM-5) criteria (American Psychiatric Association, 2022) with a diagnostic confirmation based on the Mini International Neuropsychiatric Interview for Children and Adolescents (MINI-KID) (Sheehan et al., 2010), a baseline rating of ≥4 (moderate impairment) on the Clinical Global Impression of Severity scale modified to address ADHD (CGI-S–ADHD) (Guy, 1976), and a baseline minimum symptoms total raw score of ≥28 on the ADHD Rating Scale Version 5 (ADHD-RS-5) (DuPaul et al., 2016).

Written informed consent was obtained from all parents/legal guardians prior to screening, and assent from all participants ≥7 years of age was obtained prior to any trial procedures.

Interventions

In both trials, participants aged 6–17 years were randomly assigned (1:1:1 ratio) to receive either low-dose extended-release centanafadine, high-dose extended-release centanafadine, or placebo, without an initial titration. Randomization was stratified by study site (both trials) and age group (in the trial in children, where 4- to 5-year-olds were a separate cohort [data to be presented separately]). Low-dose centanafadine did not meet the primary endpoint in either trial; thus, low-dose centanafadine has been excluded from this presentation of secondary and/or other efficacy endpoints. Dosing for children was weight-informed; those weighing 20 to <35 kg received 164.4 mg, those weighing 35–50 kg received 246.6 mg, and children >50 kg received 328.8 mg. Data derived from these three groups (high-dose centanafadine) were pooled for analyses (excluding data from the 4- to 5-year-old cohort). High-dose centanafadine dosing for adolescents was 328.8 mg once daily.

Outcomes

The primary efficacy endpoint for each trial (change from baseline in the ADHD-RS-5 symptoms’ total raw score at Week 6) and key secondary and other efficacy endpoints (change from baseline in the CGI-S–ADHD at Week 6 and change from baseline in the Conners 3–Parent Short Content Scale T-scores at Week 6) have been previously reported (Ward et al., 2025a, 2025b).

For these analyses, the key data variable was the change from baseline in the Connors 3–Parent Short Content Scale Executive Functioning T-scores and in the Connors 3–Parent Short and Conners 3–Self-Report Short Content Scale Learning Problems T-scores, over Weeks 1–6, which were analyzed using a mixed-effect model for repeated measures (MMRM). Line items used to assess these categories can be found in Figure 1.

FIG. 1.

Items used to assess executive functioning and learning problems in children and adolescents. The Conners 3–Self-Report Short does not assess executive functioning.

Additional outcome measures were clinically meaningful within-patient change in the Connors 3–Parent Short Content Scale T-scores (both post hoc analyses) and findings from the respective exit survey conducted for each trial relating to completing tasks at home, completing work at school, and the ability to learn. An anchor-based approach was used to determine the thresholds for clinically meaningful within-patient change. An anchor-based approach is a method used to interpret the clinical meaningfulness of changes in an outcome measure by relating those changes to an external criterion, or “anchor,” that is itself clinically interpretable. The anchor serves as a reference point to define what magnitude of change corresponds to a meaningful improvement or deterioration. This approach helps translate statistically significant changes into changes that are relevant and understandable in real-world clinical terms. In this study, a mean change score analysis on the Conners Content Scale T-scores from baseline to Week 6 was conducted by comparing two anchors—the CGI-S–ADHD and Patient Global Impression of Severity—where a two-point improvement was considered a meaningful improvement at Week 6. Spearman correlations between change on each of the Conners Content Scales were calculated for children and adolescents (data on file, Otsuka). For Executive Functioning, this was defined as a ≥13-point improvement in the Conners 3–Parent Short because at the population level, this would correspond to a two-point, clinically meaningful change in CGI-S–ADHD score (data on file, Otsuka). For Learning Problems, it was defined as a ≥9-point improvement in the Conners 3–Parent Short (data on file, Otsuka).

An entry survey (baseline) and an exit survey (Week 6 or trial completion), consisting of questions pertaining to unmet needs, treatment history, expectations, and outcomes of interest, were administered to parents/caregivers of all participants and to adolescent participants. The exit survey question, “How did the study medication affect each of the following areas of your/your child’s life? (Please select one response for each row),” had responses including “much worse,” “somewhat worse,” “no change,” “somewhat better,” and “much better.”

Safety was previously reported and included monitoring using standard assessments, including the frequency and severity of adverse events (AEs), and the results of laboratory tests, physical examination, vital signs measurement, and electrocardiograms (Ward et al., 2025a, 2025b).

Statistical analyses

The calculations of the sample sizes to detect the primary efficacy endpoint have been described (Ward et al., 2025a, 2025b). In brief, to control the overall experiment-wise type I error at the 0.05 level, the analysis began with a global test of the average effect of both centanafadine groups (high-dose and low-dose). If the global test was statistically significant, then individual comparisons of centanafadine low-dose or high-dose to placebo were performed in a parallel way. After both the global test and the two individual comparisons were significant for the primary efficacy endpoint at an alpha level of 0.05, then the fixed-sequence testing approach was planned for key secondary efficacy endpoints (Ward et al., 2025a, 2025b).

Data for this analysis were obtained from the participants who were assigned to high-dose centanafadine (children, n = 162; adolescents, n = 155) and placebo (children, n = 164; adolescents, n = 149) arms. For individual Conners scales, the analysis was performed on randomly assigned participants who had baseline and at least one postbaseline assessment for that scale. For the observed cases dataset (i.e., with no imputation for missing data), efficacy endpoints for centanafadine versus placebo were analyzed using an MMRM with an unstructured covariance, with trial site, treatment group, visit, and treatment group-by-visit interaction as factors and baseline-by-visit interaction as a covariate. Least squares mean and standard error for the change from baseline in Conners 3–PS Executive Functioning T-Scores and Learning Problems T-Score were estimated for each visit. The percentage of participants in each arm who achieved clinically meaningful within-patient change over time in Conners 3–Parent Short Executive Functioning T-Scores and Conners 3–Parent Short Learning Problems T-Scores was analyzed via a Cochran–Mantel–Haenszel test on the last observation carried forward analysis set. Survey data were reported using summary statistics, with values grouped for those who reported answers of “somewhat better” and “much better.”

p-Values were exploratory and were not controlled for multiplicity. The effect size, using Cohen’s d method based on MMRM estimates, was reported (Cortina and Nouri, 2000).

Results

Participants

Overall, 76.5% (367/480) of children and 80.8% (371/459) of adolescents completed their respective studies. For children, 200/480 participants (41.7%) were female with a mean age of 9.2 years; for adolescents, 187/459 participants (40.7%) were female with a mean age of 14.7 years. Baseline demographic and clinical characteristics are reported (Table 1). In children, the mean baseline (standard deviation) ADHD-RS-5 total scores were 42.9 (6.6) (centanafadine) and 43.2 (6.8) (placebo), and the mean CGI-S–ADHD scores were 4.8 (0.7) and 4.9 (0.7), respectively, indicating in aggregate that participants were markedly ill. In adolescents, the mean baseline ADHD-RS-5 total scores were 37.8 (6.4) (centanafadine) and 37.0 (5.7) (placebo), and mean CGI-S–ADHD scores were 4.5 (0.6) and 4.4 (0.5), respectively, indicating that participants were moderately to markedly ill.

Table 1.

Baseline Demographics and Clinical Characteristics

	Children		Adolescents
	Centanafadine n = 162	Placebo n = 164	Centanafadine n = 155	Placebo n = 149
Age (years), mean (SD)	9.1 (2.0)	9.2 (2.1)	14.7 (1.4)	14.7 (1.3)
Range	6–12	6–12	13–17	13–17
Sex at birth, n (%)
Female	75 (46.3)	58 (35.4)	56 (36.1)	62 (41.6)
Male	87 (53.7)	106 (64.6)	99 (63.9)	87 (58.4)
Weight (kg), mean (SD)	40.8 (15.2)	40.8 (16.0)	69.0 (19.7)	67.5 (16.9)
BMI (kg/m²), mean (SD)	20.3 (4.9)	19.8 (4.5)	24.3 (5.6)	24.2 (5.3)
Race, n (%)
American Indian or Alaskan Native	1 (0.6)	0	1 (0.6)	3 (2.0)
Asian	2 (1.2)	2 (1.2)	1 (0.6)	1 (0.7)
Black or African American	47 (29.0)	44 (26.8)	36 (23.2)	35 (23.5)
Native Hawaiian or other Pacific islander	0	2 (1.2)	1 (0.6)	0
Other	7 (4.3)	11 (6.7)	6 (3.9)	3 (2.0)
White	105 (64.8)	105 (64.0)	110 (71.0)	107 (71.8)
Conners 3–Parent Short, mean (SD)
Learning problems	70.5 (13.2)^a	70.2 (13.8)^b	66.3 (11.9)^c	66.1 (12.9)^d
Executive function	78.1 (11.5)^a	78.8 (11.0)^b	74.4 (10.4)^c	74.5 (10.5)^d
Conners 3–Self-Report Short, mean (SD)^g
Learning problems	—	—	63.2 (14.8)^e	63.0 (11.7)^f

Data are from the randomized analysis set.

Mean (SD) was calculated for ^a n = 156, ^b n = 161, ^c n = 149, ^d n = 146, ^e n = 150, and ^f n = 146.

Conners 3–Self-Report Short Executive Function Content Scale was not performed for adolescents.

BMI, body mass index; SD, standard deviation.

Efficacy

At Week 6, there were clinically significant improvements in executive functioning on the Conners 3–Parent Short Executive Functioning content scale with centanafadine compared to placebo in both children (Fig. 2A; T-score least squares mean change from baseline [standard error]: −6.8 [1.1] vs. −11.3 [1.1], p = 0.0026) and adolescents (Fig. 2B; −8.1 [1.0] vs. −13.0 [1.0], p = 0.0003). Between-arm differences were significant from Week 1 in children and from Week 2 in adolescents. At Week 6, the effect size was 0.35 for children and 0.44 for adolescents on the Conners 3–Parent Short Executive Functioning Content Scale.

FIG. 2.

Change from baseline in Conners Executive Functioning Content Scale T-scores for (A) children and (B) adolescents. *p ≤ 0.05, **p ≤ 0.01, and ***p ≤ 0.001. Mean change was derived from a mixed-effect model for repeated measures. Error bars are the standard error of the means. Data are from the modified intent-to-treat population who had a baseline value and at least 1 valid postrandomization efficacy evaluation for ADHD-RS-5 symptoms total raw score in the double-blind treatment period. ADHD, attention-deficit/hyperactivity disorder; ADHD-RS-5, ADHD Rating Scale Version 5; CTN, centanafadine; LSM, least squares mean.

Similarly, at Week 6, mean change from baseline in Conners–3 Learning Problems Content Scale T-scores was also greater in the centanafadine arms compared with placebo in both children (−8.2 [1.0] vs. −2.8 [0.9], p < 0.0001) and adolescents (−8.0 [0.9] vs. −3.1 [0.9], p < 0.0001) on the Conners 3–Parent Short content scale and in adolescents on the Conners 3–Self-Report Short scale (−6.1 [0.9] vs. −2.5 [0.9], p = 0.0023) (Fig. 3). Between-arm differences on the Conners 3–Parent Short in children were significant from Week 1. In adolescents, differences were significant from Week 2 on the Conners 3–Parent Short and from Week 4 on the Conners 3–Self-Report Short. At Week 6, the effect size was 0.50 for children and 0.48 for adolescents on the Conners 3–Parent Short Learning Problems Content Scale. A similar trend was observed for adolescents on the Conners 3–Self-Report Short Learning Problems Content Scale.

FIG. 3.

Change from baseline in Conners Learning Problems Content Scale T-scores for (A) children (Parent Short), (B) adolescents (Parent Short), and (C) adolescents (Self-Report Short). *p ≤ 0.05, **p ≤ 0.01, and ***p ≤ 0.001. Mean change was derived from a mixed-effect model for repeated measures. Error bars are the standard error of the means. Data are from the modified intent-to-treat population who had a baseline value and at least 1 valid postrandomization efficacy evaluation for ADHD-RS-5 symptoms’ total raw score in the double-blind treatment period. ADHD, attention-deficit/hyperactivity disorder; ADHD-RS-5, ADHD Rating Scale Version 5; CTN, centanafadine; LSM, least squares mean.

The percentages of participants with clinically meaningful within-patient changes up to Week 6 relative to baseline in the Conners 3–Parent Short Executive Functioning and Learning Problems Content Scale T-scores were significantly greater with centanafadine versus placebo in both children (Fig. 4A) and adolescents (Fig. 4B). Compared to placebo, children treated with centanafadine had a 50% higher likelihood (39.5% vs. 26.2%), and adolescents had an 88% higher likelihood (48.2% vs. 26.2%), of experiencing clinically meaningful within-patient change in executive functioning. Similarly, children treated with centanafadine had a 79% higher likelihood (38.8% vs. 21.5%), and adolescents had an 81% higher likelihood (40.4% vs. 24.1%), of experiencing clinically meaningful within-patient change in learning problems when compared to placebo.

FIG. 4.

Clinically meaningful within-patient changes in Conners 3–Parent Short Executive Functioning and Learning Problems Content Scale T-Scores for (A) children and (B) adolescents. *p ≤ 0.05, **p ≤ 0.01, and ***p ≤ 0.001. ^a≥13-point improvement in Conners 3–Parent Short. ^b≥9-point improvement in Conners 3–Parent Short. An anchor-based approach was used to determine the thresholds for clinically meaningful within-patient change, where a mean change score analysis on the Conners Content Scale T-scores from baseline to Week 6 was conducted by comparing to two anchors—the CGI-S–ADHD and PGI-S–ADHD. Evaluated using the Cochran–Mantel–Haenszel general association test, controlling for trial site, in a last observation carried forward analysis. CGI-S–ADHD, Clinical Global Impression of Severity scale modified to address attention-deficit/hyperactivity disorder; CI, confidence interval; CTN, centanafadine; PGI-S–ADHD, Patient Global Impression—Severity—ADHD version; RR, relative risk.

The proportions of participants who reported that they were “somewhat better” or “much better” during the exit survey in terms of completing tasks at home, completing work at school, and ability to learn were all numerically greater with centanafadine compared with placebo (Fig. 5). Data for the other outcomes (“no change,” “somewhat worse,” and “much worse”) are reported in Supplementary Table S1.

FIG. 5.

Impact of centanafadine on completing tasks at home, completing work at school, and ability to learn in (A) children (parent/caregiver report), (B) adolescents (parent/caregiver report), and (C) adolescents (self-report). Survey Question: How did the study medication affect each of the following areas of your/your child’s life? (Please select one response for each row.) Reponses included “much worse,” “somewhat worse,” “no change,” “somewhat better,” and “much better.” Values here reflect those who selected either “somewhat better” or “much better.” CTN, centanafadine.

Discussion

We have previously shown that centanafadine is efficacious and well tolerated in the treatment of ADHD in children and adolescents (Ward et al., 2025a, 2025b). The current analyses show that centanafadine clinically and significantly improved executive functioning and learning problems in the majority of treated children and adolescents with ADHD. Of note, these analyses were conducted in the overall study population rather than a cohort enriched for high executive dysfunction at baseline, yet significant improvements were still observed. Notably, both caregivers (of children and adolescents) and adolescents themselves had greater perceptions of symptom improvement with centanafadine. These data suggest that centanafadine is not only effective for treating core symptoms of ADHD but also symptoms of executive dysfunction in children and adolescents with ADHD.

Executive dysfunction can lead to difficulties planning, organizing, and completing tasks (Kosheleff et al., 2023), which can impact learning, cause deficits in daily school activities such as reading comprehension (Avramovich and Yeari, 2024; Miranda-Casas et al., 2010) and arithmetic (Kaskens et al., 2022; Viterbori et al., 2017), and lead to restricted educational attainment and subsequent employability (Varrasi et al., 2022). Stimulant and nonstimulant medications have been shown to variably improve components of executive functioning in children and adolescents with ADHD (Elliott et al., 2020; Isfandnia et al., 2024; Rosenau et al., 2021). However, stimulants have been found to have small-to-moderate effect sizes when treating executive functioning compared to core ADHD symptoms, resulting in residual impairment (Biederman et al., 2011; Isfandnia et al., 2024; Rubia et al., 2014). Moreover, greater baseline executive dysfunction is associated with a less robust ADHD response to stimulant treatment (Findling et al., 2013).

Pharmacological research on ADHD has primarily highlighted the roles of norepinephrine and dopamine in the neurobiology of ADHD (Faraone, 2009, 2018). Emerging evidence has shown that serotonergic activity interacts with dopaminergic and noradrenergic pathways, suggesting that serotonin may play a key role in the pathophysiology of ADHD (da Silva et al., 2023; Hou et al., 2018; Pourhamzeh et al., 2022). These monoamines have been found to underlie not only executive functioning but also other associated features and common comorbidities of ADHD, including sleep, memory, mood regulation, anxiety, and depression (Biederman and Spencer, 1999; Lucki, 1998; MacDonald et al., 2024; Pourhamzeh et al., 2022). Centanafadine has a broader mechanism of action compared to most existing ADHD treatments (Bymaster et al., 2012), which may explain the robust reduction in executive dysfunction symptoms. While the full impact of its mechanism of action has yet to be elucidated, it appears that centanafadine manifests therapeutic effects on executive functioning in children and adolescents with ADHD.

Limitations of this study relate to the way that executive functioning was measured in the two trials. Executive functioning is broad, and the Conners 3 assessments were able to capture only some components. Assessment of executive functioning can be done utilizing both clinical and neuropsychological scales, with the two providing different insights into a patient’s overall functional status (Ward et al., 2025b). In the current study, we employed only clinical measurements. Clinical outcome measures, such as the Conners 3–Parent Short and –Self-Report Short, often involve self-report or observer-report questionnaires that capture everyday executive functioning in real-world contexts (Conners, 2008) In contrast, neuropsychological assessments typically involve more objective, performance-based tasks that isolate specific cognitive processes under standardized conditions (Wilens et al., 2024). Combining both clinical and neuropsychological measures may offer a more holistic evaluation of executive functioning (Wilens et al., 2024). This article reports outcomes for the total trial populations rather than for a subgroup enriched for high executive dysfunction at baseline. As a result, participants with less executive dysfunction were included in analyses of change, which may have limited sensitivity to detect treatment-related improvements and introduced potential ceiling effects. Of interest, centanafadine-treated participants showed improvements using both the Conners 3–Parent Short and –Self-Report Short scales and an exit survey questionnaire in our studies, which adds support to our overall conclusions. Moreover, the samples evaluated were derived from carefully screened participants entering clinical trials and may not generalize to youth in general practice. Another limitation of this study is the absence of a formal assessment of blinding fidelity. Participants or caregivers were not asked to guess treatment allocation, nor were postrandomization expectations of response measured. As a result, it is not possible to determine whether blinding was fully maintained or whether AEs in the active treatment groups may have contributed to unblinding and influenced subjective outcome ratings. Finally, because this study did not include head-to-head comparisons between centanafadine and currently approved medications or stimulants with respect to executive functioning outcomes, the findings can only be interpreted as suggesting a potential advantage of centanafadine for executive dysfunction. Confirmation of this potential benefit will require additional studies specifically designed to directly compare centanafadine with established treatments. Furthermore, whether observed improvements in putative measures of executive functioning translate into meaningful real-world outcomes, such as academic performance, remains speculative and warrants further investigation.

Conclusions

Despite the limitations of these analyses, in addition to its impact on the core symptoms of ADHD (Ward et al., 2025a, 2025b), centanafadine appears effective in improving executive functioning, learning problems, and impact on daily tasks in children and adolescents. These data suggest that centanafadine can also benefit common associated features of ADHD and improve daily functioning, such as completing tasks at home and school. Future studies to further characterize the impact of executive function in individuals receiving centanafadine would be beneficial.

Clinical Significance

ADHD is defined by the core symptoms of inattention, hyperactivity, and impulsivity; however, there are also many associated features of ADHD, such as executive dysfunction or emotional dysregulation, that can cause a large amount of impairment for individuals living with ADHD. Previous studies have demonstrated that centanafadine was efficacious in treating the core symptoms of ADHD in children and adolescents. This study demonstrates that centanafadine also has a favorable efficacy profile in addressing associated features of executive functioning and learning problems in children and adolescents with ADHD in both caregiver-reported and adolescent self-reported outcome measures. Centanafadine is an NDSRI that differentiates it from other currently available therapeutic options for ADHD. These data presented here support that centanafadine could potentially be a therapeutic option to fill the current treatment gaps for executive functioning associated with ADHD.

Authors’ Contributions

Conceptualization: C.L.W., A.C.C., T.E.W., and D.O. contributed to the study concept and design. Methodology: C.L.W., A.C.C., T.E.W., and D.O. developed the methodology and study protocols. Formal analysis: N.J. performed the formal analysis of the data. Investigation: C.L.W., A.C.C., J.v.S., and T.S. conducted experiments, data collection, or investigation. Resources: T.S. provided essential resources, materials, or patient data. Data curation: T.S. curated, organized, and maintained the dataset. Writing—original draft: All authors drafted the initial article. Writing—review and editing: All authors reviewed and edited the article. Visualization: All authors prepared figures, tables, and visual representations of data. Supervision: C.L.W. and T.S. supervised the project and guided the research activities. Project administration: C.L.W. managed the overall project and coordinated contributions.

Footnotes

Acknowledgments

The authors thank Sally-Anne Mitchell, PhD, and Julia L. Jones, PhD, of The Medicine Group, LLC (New Hope, PA, USA), for providing medical writing support in accordance with Good Publication Practice guidelines. Medical writing support was funded by Otsuka Pharmaceutical Development & Commercialization, Inc., USA. Some of these data were previously presented at the American Professional Society of ADHD and Related Disorders (APSARD) 2025 annual conference (January 16–19, 2025; San Diego, CA, USA).

Data Sharing Statement

To submit inquiries related to Otsuka clinical research or to request access to individual participant data (IPD) associated with any Otsuka clinical trial, please visit . For all approved IPD access requests, Otsuka will share anonymized IPD on a remotely accessible data sharing platform.

Disclosures

C.L.W., D.O., N.J., and T.S. are all current employees of Otsuka Pharmaceutical Development & Commercialization, Inc., Rockville, Maryland, USA. A.C.C. has been a consultant for Aardvark, Alora, Arbor, Attentiv, Aytu, Corium, Ironshore, Lumos, Neos Therapeutics, Neurocentria, Noven, Otsuka, Purdue, Rhodes, Sky, Sunovion, Supernus, Tris, and Zevra Therapeutics Inc. (previously KemPharm Inc.); participated on speakers’ bureaus for Collegium (Ironshore), Supernus, Takeda, and Tris; has received research support from Aardvark, Adlon, Akili, Allergan, Arbor, Emalex, IntraCellular Therapies, Collegium (Ironshore), Lumos, Neos Therapeutics, Otsuka, Purdue, Rhodes, Servier, Sunovion, Supernus, Suven, Syndeio, Takeda, Tris, and Zevra Therapeutics Inc. (previously KemPharm Inc.); has received writing support from Arbor, Collegium (Ironshore), Neos Therapeutics, Purdue, Rhodes, Sunovion, Takeda, and Tris; and participated on advisory boards for Adlon, Akili, Arbor, Cingulate, Corium, Collegium (Ironshore), Otsuka, Purdue, Sunovion, Supernus, and Tris. J.v.S. has received consulting fees from Janssen, Otsuka, Purdue, and Takeda; has received advisory board fees, educational grants, and speaker fees from Janssen, Purdue, and Takeda; owns Johnson & Johnson stock; has received research grants from Biohaven, Emalex, GW Research Ltd., Janssen, Nuvelution, Otsuka, and Teva; and has received funding for multicenter trials as a primary investigator from Elvium, Purdue, and Takeda. T.E.W. has received grant/research support from NIH (NIDA) as a principal investigator; receives royalties and owns intellectual property with Cambridge University Press, Guilford Press, and 3D Therapeutics; is a consultant for Bay Cove Human Services, Gavin Foundation, and U.S. Minor/Major League Baseball; and is a coeditor for Elsevier Psychiatric Clinics of North America (ADHD).

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